Authors' Reply

  • Hung D Nguyen,
  • Maki Yoshihama,
  • Naoya Kenmochi
  • Published: July 28, 2006
  • DOI: 10.1371/journal.pcbi.0020083

Authors' Reply

We thank Csűrös for his thoughtful remarks about our recent paper in PLoS Computational Biology [1,2]. Although both our method [2] and that of Csűrös [3] (the latter was published when the former was under review) assume the same model of intron evolution, and the results obtained by both methods for the dataset in [4] are similar, the implementation details are quite different. Csűrös [3] used a trial-and-error procedure for guessing the number of unobserved intron sites, and used posterior probability calculation to optimize the rates of intron gain and loss. In contrast, in our method the number of unobserved intron sites and rates of intron gain and loss are inferred at the same time by maximizing a likelihood function. As pointed out by Csűrös [1], the recent result of Raible et al. in Science [5], which indicated that about two-thirds of Platynereis dumerilii introns are at the same positions as those in humans seems to support the results in [2] and [3]. It is likely that the evolution of introns in P. dumerilii was similar to that in humans, where two-thirds of the introns were already present in the last common ancestor and the remaining one-third was gained late after divergence from this ancestor.

Csűrös [1] commented that his algorithm is more efficient in terms of running time compared with ours. This comment is correct but with the stipulation that the number of observed sites be less than 2N, where N is the number of species. This condition does not hold for the dataset in [4]. In practice, however, this condition often holds for large values of N. In this case, some of the equations in our method [2] can be rewritten to yield the same time complexity as the method in [3].

Suppose that there are U observed intron sites that belong to V (VU) intron patterns. Denote ni (i = 0..V) to be the number of intron sites for pattern i. Note that n0, which is the number of unobserved intron sites, is unknown. The log-likelihood function in our method (Equation 7 in [2]) can be now rewritten as:
where pi is the expected probability of intron sites of pattern i and P = U + n0 is the total number of sites. Although the time to compute pi (Equation 5 in [2]) appears to grow exponentially with N, in fact it can be computed in linear time using the well-known “pruning” technique of Felsenstein [6]. For each pattern i and each node x we compute the two conditional likelihoods Lix(0) and Lix(1) for states 0 (intron absence) and 1 (intron presence), respectively, using the post-order tree traversal. After that we compute:
where λ is the probability of introns being present at the root node and Lir(0) and Lir(1) are the two conditional likelihoods of the root node for pattern i.

Since all the conditional likelihoods for every node are now already known, the expected counts of intron gain and loss for each intron pattern i along each branch k, gik, and lik, as well as the expected intron counts at each node h, oih, can also be computed in linear time with N using the pruning technique, but this time with the pre-order tree traversal. Finally, Equations 9 and 10 in [2] can be rewritten as:
where and are, respectively, the conditional expected counts of intron gain and loss for each branch k given the data; is the conditional expected intron count for each node h given the data; and is the expected number of sites for each intron pattern i. In this way, our algorithm also has a linear time with N and V. In fact, the code that was released together with our paper [2] has already been implemented using the pruning technique.

Csűrös [1] also commented that our Proposition 1 echoes the Pulley Principle of Felsenstein [6] for ambiguous root placement. The Pulley Principle, however, applies to only reversible Markov processes whereas our model of intron evolution is an irreversible one. Concerning our Proposition 2, Csűrös [1] is correct to comment that our method for finding the most biologically meaningful solution, which is based on the variance of intron gains and losses, is less efficient. Since our algorithm is initialized with very small rates of intron gain and loss, the algorithm almost always converges to the most biologically meaningful solution (although there is no direct proof for this). Thus, the step of finding the most biologically meaningful solution was added to make the algorithm more rigorous, and may be removed in practice.

Although the use of Equation 1 in [1] will lead to a unique solution with the method of Csűrös, the method may not always find a solution with the maximum likelihood. This happens when no solution among the 2N−2 optimal solutions (see our Proposition 2 in [2]) satisfies the condition pe(0 → 1) + pe(1 → 0) < 1 in [1] (or αk + βk < 1 in our terms, where αk and βk are the probabilities of intron gain and loss along branch k, respectively [2]). Let us consider the following example using Figure S3 in [2]. Suppose that C and D are external nodes (i.e., they present the observed data) and oC = 420 and oD = 160, where oX shows the number of introns at node X. We suppose further that P = 1,000 and that one optimal solution for node B has the following parameters: oB = 400, gy = 60, ly = 40, gz = 120, and lz = 360. In this case, αy = 0.1, βy = 0.1, αz = 0.2, βz = 0.9, and αz + βz > 1. According to our Protocol S2, the other optimal solution for node B has the following parameters: oB = 600, gy = 360, ly = 540, gz = 40, and lz = 480. In this case, αy = 0.9, βy = 0.9, αz = 0.1, βz = 0.8, and αy + βy > 1. That is, both optimal solutions for node B violate the condition αk + βk < 1, and the method of Csűrös may not find an optimal solution in this case. Therefore, to always find the most biologically meaningful solution in linear time using our method, we propose to use a new definition: the most biologically meaningful solution is the one that has the least total number of intron gains and losses. Now we can compute the most biologically meaningful solution from any arbitrary optimal solutions by using the post-order tree traversal, and for each internal node B choose the optimal solution for which the sum gy + ly + gz + lz is smaller (i.e., the case oB = 400 in the above example). The algorithm clearly has a linear time with N.

Csűrös [3] stated that it was not possible to estimate n0 (i.e., the number of unobserved intron sites) by means of likelihood. Therefore, a trial-and-error procedure, which basically tries all possible values for n0, was used. Our method, however, suggests that n0 can be optimized by means of likelihood. The problem may be that the method in [3] attempts to maximize a log-likelihood function similar to the one in Equation 1 (in this reply) but without the last two terms. When n0 is invariant, these two terms are constant and can be omitted. However, they cannot be omitted when inferring intron evolution (where n0 is unknown) if we want to optimize n0 by means of likelihood. One advantage of using our log-likelihood function is that we can use conventional methods (such as the Brent algorithm) for optimizing n0, which are more efficient than the trial-and-error method employed in [3]. Another advantage of using our function is that different trees can be compared on the basis of their likelihoods [2]. The method in [3] does not allow such a comparison.

Our intent when comparing the number of possible patterns with the number of intron sites in the dataset in [7] was to show that the dataset may be insufficiently large for a valid inference, i.e., other methods such as maximum parsimony may perform better in this case. It was not our intent to claim that the sample data must grow proportionally with the number of possible patterns for a statistical inference to be valid. Our recent simulations (unpublished data) seem to support our speculations about the dataset in [7].


  1. 1. Csűrös M (2006) On the estimation of intron evolution. PLoS Comput Biol 2: e84.. DOI: 10.1371/journal.pcbi.0020084.
  2. 2. Nguyen HD, Yoshihama M, Kenmochi N (2005) New maximum likelihood estimators for eukaryotic intron evolution. PLoS Comput Biol 1: e79.. DOI: 10.1371/journal.pcbi.0010079.
  3. 3. Csűrös M (2005) Likely scenarios of intron evolution. In: McLysaght A, Huson D, editors. Proceedings of Comparative Genomics: RECOMB 2005 International Workshop. Berlin: Springer-Verlag. pp. 47–60. Lecture Notes in Bioinformatics 3678.
  4. 4. Rogozin IB, Wolf YI, Sorokin AV, Mirkin BG, Koonin EV (2003) Remarkable interkingdom conservation of intron positions and massive, lineage-specific intron loss and gain in eukaryotic evolution. Curr Biol 13: 1512–1517.
  5. 5. Raible F, Tessmar-Raible K, Osoegawa K, Wincker P, Jubin C, et al. (2005) Vertebrate-type intron-rich genes in the marine annelid Platyneeis dumerilii. Science 310: 1325–1326.
  6. 6. Felsenstein J (1981) Evolutionary trees from DNA sequences: A maximum likelihood approach. J Mol Evol 17: 368–376.
  7. 7. Qiu WG, Schisler N, Stoltzfus A (2004) The evolutionary gain of spliceosomal introns: Sequence and phase preferences. Mol Biol Evol 21: 1252–1263.